Population-sample regression in the estimation of population proportions

ABSTRACT

Accordingly, embodiments of the present invention advantageously account for item discrimination in a single-parameter logistic model used for measuring a test-taker&#39;s ability and an item&#39;s difficulty. Accounting for item discrimination improves the reliability of a test without increasing the number of test items. To account for item discrimination in a single-parameter logistic model, this invention uses the correlation between item response (correct or incorrect) and total test score or other measure of test-taker ability to obtain a Bayesian estimate of the correct-response probability (between zero and one). This correlation is a measure of item discrimination. The numerator in the formula for this correlation contains the difference between the average test score of test-takers who got the item right and the average test score of test-takers who got the item wrong.

CROSS-REFERENCE TO RELATED APPLICATIONS

This application is a Continuation-in-Part of and claims benefit toco-pending, commonly owned U.S. patent application Ser. No. 13/923,217,filed Jun. 20, 2013, which was a continuation of U.S. patent applicationSer. No. 11/845,677 filed Aug. 27, 2007 which claimed the benefit ofU.S. Provisional Patent Application No. 60/823,625 filed Aug. 25, 2006,and was a continuation of U.S. patent application Ser. No. 12/972,397filed Dec. 17, 2010, which claimed the benefit of U.S. ProvisionalPatent Application No. 61/288,156 filed Dec. 18, 2009, all of which areincorporated herein by reference.

BACKGROUND OF THE INVENTION

There are numerous methods of estimating population proportions forpolls. In 1961, James and Stein derived estimators of population meansthat are more efficient than corresponding traditional estimators byusing a linear combination of the mean of an individual sample and theoverall mean of the sample aggregated with two or more other samplesfrom possibly different populations. Being within the 0-1 interval, theweight applied to each individual sample mean is called a shrinkagecoefficient.

Commenting on the empirical Bayesian treatment of James-Stein estimatorsby Efron and Morris (1973), Stigler (1983, 1990) showed that theshrinkage coefficient was an estimator of the squared correlationcoefficient in the regression of population on sample means. Fienbergand Holland (1973) extended the empirical Bayesian treatment ofJames-Stein estimators to single-sample population proportions, with theexpected increase in efficiency.

Likewise, there are numerous methods of testing the ability of a subjectand/or the difficulty of a task. Testing methods originally tended tofocus on the total test score. Over time testing methods have developedto include a focus on individual responses.

As the focus on total test scores in classical test theory shifted toindividual item responses in modern test theory, the models underlyingthe theories changed correspondingly from measurement and estimation toprobabilistic models. In the measurement model of classical test theory,an observed test score (X) differs from a true test score (T) by error(E): X=T+E. Measurement error (E) formally disappears in modern testtheory, where the concept of uncertainty expressed by responseprobability replaces the concept of imprecision expressed by measurementerror. In modern test theory, the probability of a correct response toan item is a function of an examinee's ability (θ) and the item'sdifficulty (b), as well as possibly other item parameters such as thediscrimination parameter (a) and, for multiple-choice items, theguessing-rate parameter (c): P(θ, a, b, c). While probabilistic modelsinvolving all three item parameters are popular because of their promiseof optimal fit, many test developers use the statistically simplersingle-parameter logistic model introduced by Rasch (1960):

$\begin{matrix}{{P\left( {\theta;b} \right)} = \frac{1}{1 + e^{- {({\theta - b})}}}} & (1)\end{matrix}$the graph of which is an ogive curve centered at b on the θ scale.

Prior to Rasch (1960), Birnbaum (1958) introduced logistic item responsemodels, his two-parameter version involving both the location(difficulty) parameter b and the slope (discrimination) parameter a:

$\begin{matrix}{{P\left( {{\theta;\alpha},b} \right)} = \frac{1}{1 + e^{- {\alpha{({\theta - b})}}}}} & (2)\end{matrix}$Like Equation (1), the graph of Equation (2) is an ogive on the θ scalecentered at b; but, different from Equation (1), the slope of the ogivemay vary depending on the value of the parameter a, which, in thecontext of the relationship between measurement error and responseprobability, is the focus of this disclosure.

Some test developers may consider the single-parameter model describedby Equation (1) as unnecessarily limited in its data-fitting ability incontrast to alternatively available two- or three-parameter models. Yet,studies show that the Rasch model may fit data at least as well as itsmultiple-parameter counterparts (e.g., Forsyth, Saisangjan, & Gilmer,1981). Thissen (1982), in particular, showed that the addition of theparameter a to the Rasch model may fail to improve model fitsignificantly. Because other studies may show otherwise (e.g., DeMars,2001, and Stone & Yumoto, 2004), some test developers who favor theRasch model might still wish that a single-parameter logistic modelcould accommodate differences in item discrimination, as well as itemdifficulty. At the same time, the allowance of differences in itemdiscrimination to affect the estimation of θ values may disturb othersupporters of the Rasch model because they believe it unfair to weightresponses to items differently, at least without informing test-takers.That concern, however, cannot justify counting clearly less and morediscriminating items equally in scoring, particularly when the resultsof equal and appropriate unequal weighting of item responses differsubstantially.

Different from the development here, involving estimation of itemdiscrimination from data, Verhelst and Glas (1995) introduced thediscrimination parameter a_(i) into the Rasch model as an unestimatedconstant to account for varying item discrimination. Because itexplicitly lacked the unweighted-scores property of the Rasch model,they also referred to their model simply as a single-parameter logisticmodel. Weitzman (1996) used an adjustment of x_(iq) like p_(iq) (seebelow, Detailed Description of Invention) to enable the Rasch model toaccount for guessing, but that adjustment required the assumption thatthe guessing rate was constant over items. Weitzman (2009) provides theoriginal account of the invention described here.

Generally, single-parameter models do not tend to account for itemdiscrimination, which is how well the item measures what it is supposedto measure. Single-parameter models do, however, lead to accurateequating of different test forms. Two and three parameters models tendto account for item discrimination. However, two and three parametermodels lead to inaccurate equating of different test forms. Accordingly,a need exists for improved test modeling.

SUMMARY OF THE INVENTION

The present technology may best be understood by referring to thefollowing description and accompanying drawings that are used toillustrate embodiments of the present technology directed towardapplication of regression of a true measurement on an observedmeasurement to statistical estimation of the true measurement.

In a population proportion on sample proportion regression embodiment ofthe present technology, a method for determining the results of a pollhaving a smaller than conventional margin of error for any sample sizeor a smaller than conventional sample size for any margin of error, themethod comprising statistically determining the regression of populationon sample proportions by using an unbiased estimator of the square ofthe correlation between them. Applying this method to a single sampleobtained randomly with replacement from a single population results incredibility intervals that are narrower than conventional confidenceintervals, shown to be a product of the regression of sample onpopulation proportions. Not only does the squared-correlation estimatorfunction as a shrinkage coefficient in a Bayesian context, but also thiscorrespondence is shown to apply generally to the estimation ofpopulation means as well as proportions. In a preferred embodiment,population-sample regression is used to develop correspondingfrequentist and Bayesian estimators.

In another embodiment, a method is presented for estimating populationfrom sample proportions that produces margins of error narrower for anyspecific sample size or that requires a sample size smaller for anyspecific margin of error than do previously existing methods applied tothe same data. This method applies an unbiased estimator of the squaredcorrelation between population and sample proportions to determine pointand interval estimates of population proportions in a regression contextinvolving simple random sampling with replacement. In virtually allreasonable applications, assuming a Dirichlet prior distribution, themargin of error produced by this method for a population proportion isshown to be 1.96 times the posterior standard deviation of theproportion.

In regressing a probability of a correct answer on an actual response inan embodiment of the present technology, a delta single-parameterlogistic modeling technique is described. The delta single-parameterlogistic model includes receiving a response for each item and eachtest-taker, and a total test score for each test-taker. A correlationbetween the item response and the total test score over the plurality oftest-takers is determined. Using this correlation, which measures itemdiscrimination, a Bayesian estimate of the probability of a correctresponse by each test-taker to each item is determined. The logit ofthis probability estimate is computed. The difficulty of each item isestimated as a function of the average logit over the number oftest-takers. Each test-taker's ability is also estimated as a functionof the average logit and the average difficulty over the number of testitems.

In another embodiment, substituting the latest estimated test-takerability for the total test score, the delta single-parameter logisticmodeling technique is iteratively performed to improve the estimate ofthe test-taker's ability. In yet another embodiment of the presenttechnology, the delta single-parameter logistic modeling technique iscombined with a Rasch modeling technique to improve the estimate of thetest-taker's ability and the difficulty of the task.

Accordingly, embodiments of the present invention advantageously accountfor item discrimination in a single-parameter logistic model used formeasuring a test-taker's ability and an item's difficulty. Accountingfor item discrimination improves the reliability of a test withoutincreasing the number of test items. To account for item discriminationin a single-parameter logistic model, this invention uses thecorrelation between item response (correct or incorrect) and total testscore or other measure of test-taker ability to obtain a Bayesianestimate of the correct-response probability (between zero and one).This correlation is a measure of item discrimination. The numerator inthe formula for this correlation contains the difference between theaverage test score of test-takers who got the item right and the averagetest score of test-takers who got the item wrong.

BRIEF DESCRIPTION OF THE DRAWINGS

Embodiments of the present technology are illustrated by way of exampleand not by way of limitation, in the figures of the accompanyingdrawings and in which like reference numerals refer to similar elementsand in which:

FIG. 1 illustrates the regression of the sample proportion P on thepopulation proportion n.

FIG. 2 illustrates the regression of the population proportion π on thesample proportion P.

FIG. 3 illustrates Bayesian (π-on-P) as a percentage of frequentitsterror margins in a two-option case.

FIG. 4 illustrates Bayesian (π-on-P) error margins in the two-optioncase.

FIG. 5 shows a flow diagram of a π-on-P point and interval estimation.

FIG. 6 shows credibility-interval coverage of ±1.96 standard errors inmost applications of π-on-P estimation.

FIG. 7 shows a comparison of π-on-P with other estimates in a particularexample involving three options.

FIG. 8 shows a comparison of critical with Bayesian and conventionalmargins of error.

FIG. 9 shows a comparison of samples sizes required to produce commonlyused error margins.

FIGS. 10A and 10B show a flow diagram of delta single-parameter logisticmodeling methods, in accordance with a number of embodiments of thepresent technology.

FIG. 11 shows a table comparing item difficulties and itemdiscrimination data for various modeling techniques.

FIG. 12 shows a graph comparing Rasch and delta single-parameterlogistic model validities.

DETAILED DESCRIPTION OF THE INVENTION

Reference will now be made in detail to the embodiments of the presenttechnology, examples of which are illustrated in the accompanyingdrawings. While the present technology will be described in conjunctionwith these embodiments, it will be understood that they are not intendedto limit the invention to these embodiments. On the contrary, theinvention is intended to cover alternatives, modifications andequivalents, which may be included within the scope of the invention asdefined by the appended claims. Furthermore, in the following detaileddescription of the present technology, numerous specific details are setforth in order to provide a thorough understanding of the presenttechnology. However, it is understood that the present technology may bepracticed without these specific details. In other instances, well-knownmethods, procedures, components, and circuits have not been described indetail as not to unnecessarily obscure aspects of the presenttechnology.

In this application, the use of the disjunctive is intended to includethe conjunctive. The use of definite or indefinite articles is notintended to indicate cardinality. In particular, a reference to “the”object or “a” object is intended to denote also one of a possibleplurality of such objects.

Embodiments of the present technology are directed toward simpleregression of a “true” measurement on an “observed” measurement. Thedirection of this regression (“true” on “observed”) corresponds toBayesian analysis, whereas the opposite direction corresponds toclassical (or standard) estimation in statistics. In one embodiment, thetrue measurement may be a population proportion, and the observedmeasurement may be a corresponding sample proportion. In anotherembodiment, the true measurement may be the probability of a correctresponse to an item by a particular person, and the observed measurementmay be the actual response by that person to the item. The simpleregression involves a correlation between the variables on the two sidesof the equation.

In population proportion-on-sample proportion regression embodiments ofthe present technology, an estimator of population proportions isdeveloped that is even more efficient than the Fienherg-Hollandestimator, particularly in a small sample (n<500), and demonstrates thatall shrinkage coefficients are estimators of squared correlationcoefficients in population-on-sample regression. The margins of errorfor the population proportions, provided they follow a Dirichletdistribution, are shown to be 1.96 times their standard deviations invirtually all realistic applications. These techniques relate to thetechnical field of poll-taking, particularly utilizing statistics whichencompass the estimation of population proportions from the regressionof population on sample proportions. Using an unbiased estimator of thesquare of the correlation between the population and sample proportionsin a Bayesian context produces not only point estimates of thepopulation proportions but also credibility intervals that are narrowerthan corresponding conventional confidence intervals.

Simple regression analysis in statistics is a procedure for estimatingthe linear relationship between a dependent variable and an independentvariable in a given population. The relationship for standardizedvariables is expressed as an equation for a straight line in which thecoefficient of the independent, or regressed-on, variable in theequation is determined from a sample. While the dependent variable mayvary, the regressed-on variable is fixed. This variable is the samplestatistic, mean or proportion, in population-on-sample regression. Thisis why population-on-sample regression corresponds to Bayesianestimation. The opposite is true of sample-on-population regression,which corresponds to frequentist estimation in which the populationparameter, mean or proportion, is fixed.

A point-estimation advantage of the population-on-sample regressionprocedure is that it generally avoids the problem of relative-frequencyestimates equal to zero or one by reasonably adjusting them away fromthese extreme values. This inward adjustment is the result of regressiontoward the mean, which for relative frequencies is greater than zero andless than one.

Population-on-sample regression shares the efficiency advantage ofBayesian over traditional estimation, which shows itself in a reductionof least-squares risk functions with a corresponding shortening ofconfidence intervals. In mental test theory, for example, standarderrors of estimate, used to determine confidence intervals for truescores, are shorter than standard errors of measurement, used todetermine confidence intervals for observed scores. The “shrinkagecoefficient” in this case is the squared true-observed-scorecorrelation, and the standard error of estimate is equal to the standarderror of measurement time this correlation.

The method comprising embodiments of this invention provides for bothpoint and interval estimation for arriving at usable results of a poll.The approach to each follows both frequentist and Bayesian tracks withina regression framework. Development of either the frequentist or theBayesian point estimator makes no distributional assumption about thepopulation proportions. Distributional assumptions come into play onlyin the treatment of interval estimation. Normally a regression approachwill require three or more observations to accommodate the need toestimate the slope and the intercept of the regression line from data.However, because the mean requires no estimation in the case ofsingle-sample proportions, being the reciprocal of the number of optionsor categories, the Bayesian as well as the frequentist point estimatordeveloped herein can apply to data involving only two (binomial ormultinomial) observations.

The focus here physically has been on a single sample obtained from asingle population. Conceptually, the sample may be one of many that thepopulation can produce or the population may be one of many that canproduce the sample. The first possibility underlies the frequentistapproach and the second the Bayesian approach to statistical inference.The method constituting embodiments of this invention has adopted theBayesian approach showing that it can lead through regression toconsiderably more efficient estimation of population proportions thanthe frequentist approach, especially for samples no larger than 500.

Although reference is made explicitly to proportions, the method appliesequally to other forms of expressing such results of a poll, forexample, percentage, fractions, and decimal fractions, with appropriateadjustments known by those skilled in the art.

The following steps develop the regression point and interval estimatorsin the frequentist (P-on-π) case (Step 1) and in the Bayesian (π-on-P)case (remaining steps).

The Regression of P on π

The frequentist approach to point estimation via regression correspondsto the traditional estimation procedure in which, for a sample of sizen, nP (an integer) has a binomial of a multinomial distribution withE_(t)(P_(kt))=π_(k) for each option k of a total of K options, tindexing the sample. The regression expressing this or the Bayesianapproach involves, for a single sample t, the mean P and μ_(π) overoptions

$\left( {\mu_{r} = {\overset{\_}{P} = {\left( \frac{1}{K} \right){\sum\limits_{k = 1}^{\; K}\; P_{kt}}}}} \right),$the standard deviations S_(P) and σπ over options

$\left( {S_{P} = \sqrt{\left( {1\text{/}K} \right){\sum\limits_{k = 1}^{K}\;\left( {P_{kt} - \mu_{P}} \right)^{2}}}} \right),$and the correlation coefficient ρ_(πP) over options. To assure thatE_(t)(P_(kt))=π_(k), the slope coefficient in the regression

$\begin{matrix}{P_{kt} = {{\left( \frac{S_{P}}{\sigma_{\pi}} \right)\rho_{\pi\; P}\pi_{k}} - {\left( \frac{S_{P}}{\sigma_{\pi}} \right)\rho_{\pi\; P}\mu_{\pi}} + \overset{\_}{P} + \epsilon_{kt}}} & (3)\end{matrix}$must be equal to one, the population and sample means μ_(π) and P beingequal to 1/K, so that P_(kt)=π_(k)+∈_(kt), where ∈_(kt) denotes samplingerror. Since E_(t)(∈_(kt))=0, E_(t)(P_(kt))=π_(k), as in the traditionalbinomial or multinomial estimation procedure. The regression implicationof E_(t)(P_(kt))=π_(k) then is that the correlation ρ_(πP) between π_(k)and P_(kt) must be equal to the ratio of their standard deviations,σ_(π) and σ_(P):

$\begin{matrix}{\rho_{\pi\; P} = \frac{\sigma_{\pi}}{S_{P}}} & (4)\end{matrix}$

This result resembles a basic result of classical mental test theory(Gulliksen, 1950) in which π represents a true and P an observed score.The next section will use this result to obtain an estimator of ρ_(πP)².

Step Two—An Estimator of ρ_(πp) ²

In this as in the previous section, for each option k, π_(k) as theregressed-on variable is assumed to be fixed while P_(kt) can vary oversamples (t=1, 2, . . . ). Ordinarily the exact value of ρ_(πP) ² isunknown. Because P_(kt) is a proportion, however, σ_(π) ² is expressedin terms of σ_(P) ², the expected value of S_(P) ², for substitutioninto Equation (4) to yield an estimator of ρ_(πP) ² in which σ_(P) ² isreplaced by S_(P) ²:

$\begin{matrix}{{S_{P}}^{2} = {\frac{1}{K}{\sum\limits_{k = 1}^{K}\;\left( {P_{kt} - \mu_{P}} \right)^{2}}}} & (5)\end{matrix}$where μ_(P) is the mean of the K values of P_(kt) in sample t. Withoutfurther assumptions or conditions, the following derivation leads to thesample estimator of ρ_(πP) ² in Equation 15.

In Equation 5, μ_(P), equal to 1/K, is the population as well as thesample mean proportion so that S_(P) ² with K rather than (K−1) in thedenominator, is an unbiased estimator of σ_(P) ²:

$\begin{matrix}{{\sigma_{P}}^{2} = {E_{t}\left( {\left( \frac{1}{K} \right){\sum\limits_{k = 1}^{K}\;{\left( {{Pkt} - \mu_{P}} \right)2}}} \right)}} & (6)\end{matrix}$If π_(k)=μ_(π)+δ_(k), where

${{\left( \frac{1}{K} \right){\sum\limits_{k = 1}^{K}\;\delta_{k}}} = 0},$then

${\sigma_{n}}^{2} = {\left( \frac{1}{K} \right){\sum\limits_{K = 1}^{k}\;{{\delta_{K}}^{2}.}}}$As noted in the preceding section, P_(kt)=π_(k)+∈_(kt). The expectedvalues of ∈_(kt) and S_(k)∈_(kt) (equal to δ_(k) times the expectedvalue of ∈_(kt)) are equal to zero. Substitution first of π_(k)+∈_(kt)for P_(kt) and then of μ_(π)+δ_(k) for π_(k) in Equation 6 thus, withμ_(π)=μ_(P), leads to

$\begin{matrix}{\sigma_{P}^{2} = {\sigma_{\pi}^{2} + {\left( \frac{1}{K} \right){\sum\limits_{k = 1}^{K}\;\sigma_{\epsilon\; k}^{2}}}}} & (7)\end{matrix}$where, for each option k, σ_(∈k) ² is the sampling variance

$\begin{matrix}{\sigma_{\epsilon\; k}^{2} = {\left( \frac{1}{n} \right){\pi_{k}\left( {1 - \pi_{k}} \right)}}} & (8)\end{matrix}$

Substitution of μ_(π)+δ_(k) for π_(k) in the computation of the mean ofEquation 8 over the K values of σ_(∈k) ² produces

$\begin{matrix}{{\frac{1}{K}{\sum\limits_{k = 1}^{K}\;{\sigma_{\epsilon\; k}}^{2}}} = {\frac{\mu_{k}\left( {1 - \mu_{k}} \right)}{n} - \frac{{\sigma_{\pi}}^{2}}{\pi}}} & (9)\end{matrix}$since

${{\sum\limits_{k = 1}^{K}\;{\mu_{\pi}\delta_{k}}} = {{\mu_{\pi}{\sum\limits_{k = 1}^{K}\;\delta_{k}}} = {{0\mspace{14mu}{and}\mspace{14mu}{\sum\limits_{k = 1}^{K}\;{\delta_{k}}^{2}}} = {K\;{\sigma_{\pi}}^{2}}}}},$Equation 7 thus becomes

$\begin{matrix}{{\sigma_{P}}^{2} = {{\sigma_{\pi}}^{2} + \frac{\mu_{\pi}\left( {1 - \mu_{\pi}} \right)}{n} - \frac{{\sigma_{\pi}}^{2}}{n}}} & (10)\end{matrix}$or, with

$\frac{1}{K}$for μ_(π),

$\begin{matrix}{{\sigma_{P}}^{2} = {{\sigma_{\pi}}^{2} + \frac{K - 1}{n\; K^{2}} - \frac{{\sigma_{\pi}}^{2}}{n}}} & (11)\end{matrix}$Solution of Equation 11 for σ_(π) ² finally yields the expression ofσ_(π) ² in terms of σ_(P) ²:

$\begin{matrix}{\mspace{20mu}{\sigma_{\pi}^{2} = {{\left( \frac{n}{n - 1} \right)\sigma_{P}^{2}} - {\left( \frac{\left. {K - 1} \right)}{n - 1} \right)\left( \frac{1}{K} \right)^{2}}}}} & (12)\end{matrix}$The formula for ρ_(πP) ² is thus

$\begin{matrix}{\rho_{\pi\; P}^{2} = \frac{{\left( \frac{n}{n - 1} \right)\sigma_{P}^{2}} - {\left( \frac{\left. {K - 1} \right)}{n - 1} \right)\left( \frac{1}{K} \right)^{2}}}{S_{P}^{2}}} & (13)\end{matrix}$so that the estimator of ρ_(πP) ² is

$\begin{matrix}{{\hat{\rho}}_{\pi\; P}^{2} = \frac{{\left( \frac{n}{n - 1} \right)S_{P}^{2}} - {\left( \frac{\left. {K - 1} \right)}{n - 1} \right)\left( \frac{1}{K} \right)^{2}}}{S_{P}^{2}}} & (14)\end{matrix}$or, since

${{S_{P}}^{2} = {{\left( \frac{1}{K} \right){\sum\limits_{k = 1}^{K}\;{P_{k}}^{2}}} - \left( \frac{1}{K} \right)^{2}}},$

$\begin{matrix}{{{\overset{.}{\rho}}_{\pi\; P}}^{2} = {1 - \frac{K\left( {1 - {\sum\limits_{k = 1}^{K}\;{P_{kt}}^{2}}} \right)}{\left( {n - 1} \right)\left( {{K{\sum\limits_{k = 1}^{K}\;{P_{kt}}^{2}}} - 1} \right)}}} & (15)\end{matrix}$where

$\sum\limits_{k = 1}^{K}\;{P_{kt}}^{2}$is the sum of the squares of the K proportions computed from the samplet of size n. Equation 14 shows that {circumflex over (ρ)}_(πP) ² is notonly an increasing function of S_(P) ², K, and n but also an unbiasedestimator of ρ_(πP) ², the S_(P) ² in the denominator being fixed.Step Three—Point Estimation: Estimation of π for P

The regressions underlying the developments in this and the precedingsection are opposite in direction. Both involve P and π. The developmentin the preceding section considered π as fixed and P as variable. In thedevelopment here, however, the reverse is true: P is fixed, and π isvariable. In this development, the fixed-P_(k) sample (k=1, 2, . . . ,K) comes from a single population, which is one of any number ofpossible populations, with their correspondingly different π_(k) values.Although the expected value of variable P_(k) is π_(k), the expectedvalue of a variable π_(k) is not P_(k), but a value {circumflex over(π)}_(k) somewhere between P_(k) and 1/K. Whereas the direction ofregression assumed in the proceeding section worked for the developmentof a formula for {circumflex over (ρ)}_(πP) ², the regression directiontaken here is particularly appropriate for the estimation of an unknownπ_(k), assumed variable, from a known P_(k), assumed fixed for eachoption k:

$\begin{matrix}{{\hat{\pi}}_{k} = {{{\rho_{\pi\; P}\left( \frac{\sigma_{P}}{S_{P}} \right)}P_{k}} - {{\rho_{\pi\; P}\left( \frac{\sigma_{P}}{S_{P}} \right)}\overset{\_}{P}} + µ_{\pi}}} & (16)\end{matrix}$where {circumflex over (π)}_(k) is the regression estimate of π_(k),ρ_(πP) is the correlation between P_(k) and π_(k) for the populationsampled from, and μ_(P) and μ_(π) are the means and S_(P) and σ_(π) arethe standard deviations over options of P_(k) and π_(k), respectively.Since ρ_(πP) ²=σ_(π) ²/S_(P) ², this equation simplifies to

$\begin{matrix}{{\hat{\pi}}_{k} = {{\rho_{\pi\; P}^{2}P_{k}} - {\rho_{\pi\; P}^{2}\overset{\_}{P}} + \mu_{\pi}}} & (17)\end{matrix}$where μ_(π) is equal to P so that, with both μ_(π) and P denoted by μ,

$\begin{matrix}{{\hat{\pi}}_{k} = {{\rho_{\pi\; P}^{2}P_{k}} + {\left( {1 - \rho_{\pi\; P}^{2}} \right)\mu}}} & (18)\end{matrix}$or, since μ=1/K,

$\begin{matrix}{{\hat{\pi}}_{k} = {{\rho_{\pi\; P}{{}_{\;}^{}{}_{}^{\;}}} + \frac{1 - \rho_{\pi\; P}^{2}}{K}}} & (19)\end{matrix}$Estimation of the population proportion π_(k) corresponding to theobserved proportion P_(k) thus requires knowledge only of ρ_(πP) ². Ifρ_(πP) ²=1, {circumflex over (π)}_(k)=P_(k); if ρ_(πP) ²=0, {circumflexover (π)}_(k)=1/K. Generally, in practice, f_(k) will be somewherebetween P_(k) and 1/K.

Since P_(k) is assumed fixed, substitution of {circumflex over (ρ)}_(πP)² for ρ_(πP) ² in Equation 19 yields an estimate of π_(k) that is notsubject to sampling variation.

FIG. 1 illustrates the regression of P on π and FIG. 2 the regression ofn on P. In both figures, the population proportions are fictitious sincetheir actual values are unknown. Knowledge of these values isunnecessary because the only requirements for estimation are the sampleproportions in FIG. 1 and the regression line in FIG. 2. The verticallines define 95% confidence intervals. Based on n=100, the value of{circumflex over (ρ)}_(πP) ² is 0.88, the slope of the regression linein FIG. 2. In addition to their different slopes, the two regressionlines notably have different intercepts: 0 in FIG. 1 and 0.03 in FIG. 2.Except when ρ_(πP) ²=1, π-on-P regression produces population-proportionestimates that are greater than zero and less than one.

Two examples provide data to illustrate the π-on-P regression procedure.The first, cited by Tull and Hawkins (1993, pp. 745-746) in the spiritof R. A. Fisher's classic tea-testing illustration of the Studentt-test, was a Carnation taste test comparing Coffee-mate to real cream.Of 285 participants who claimed to be able to distinguish between twocups of coffee presented them, one containing Coffee-mate and the othercontaining cream, 153 were correct and 132 were incorrect, thecorresponding proportions being 0.54 and 0.46. With {circumflex over(ρ)}_(πP) ²=0.42, the Bayesian-estimated proportions were 0.52 and 0.48,respectively. These proportions more strongly than their frequentistcounterparts support the conclusion that people could not tell thedifference between Coffee-mate and real cream.

In the second example, a large school district tested three differenttextbooks for first-year high school algebra. The first was the bookused for the past several years; the second and third were new bookscontaining questions taken from recent versions of a statewidemathematics examination. The question format differed in these books,being open-ended in the second book and multiple-choice (as in thestatewide examination) in the third book. Two hundred students indifferent classes used each book. Of the 450 students who passed thestatewide examination, 130 had used the first book, 158 the second book,and 162 the third book. The corresponding proportions were 0.29, 0.35,and 0.36, respectively. Substituting these proportions, which sum to1.00, in Equation 13, together with K=3 and n=450 yields 0.48 for thevalue of {circumflex over (ρ)}_(πP) ², and using this value for{circumflex over (ρ)}_(πP) ² in Equation 19 yields {circumflex over(π)}_(k) values of 0.31, 0.34 and 0.35 for the three books,respectively. The first and third values notably differ (by 0.02 each)from their uncorrected counterparts while the second, being closer tothe mean of 0.33, shows a difference of only 0.01, to two decimalplaces. If the books had been equally effective, the expectedproportions within the passing group would be equal to 0.33 for studentsusing all three books.

Step 4A—Interval Estimation: Two-Option Case

Reported survey results often include half the size of the 95%confidence interval as the so called “margin of error.” For K=2, theprocedures developed in this embodiment of the invention involveintervals different from the conventional ones. The confidence orcredibility intervals appropriate for the K=2 procedures developed hereare functions of the standard error of measurement, applicable to theregression of P on π, or the standard error of estimate, applicable tothe regression of π on P (Kelley, 1923, 1927). Both standard errorsinvolve the assumption of homoscedasticity: Values of the standarderrors of measurement are equal for all values of π_(k), and values ofthe standard error of estimate are equal for all values of P_(k) (k=1,2, 3 . . . , K). In the case of proportions, as opposed to means ofmulti-valued variables, this assumption makes sense only when K=2.

Though of less practical value, the standard error of measurement,σ_(P-π)=S_(P)√{square root over (1−ρ_(πP) ²)}, produces confidenceintervals directly comparable to the conventional ones. Estimates ofσ_(P-π) are obtainable by using {circumflex over (ρ)}_(πP) ² for ρ_(πP)² in the formula for σ_(P-π).

The Carnation data provide an example. The two observed proportions,0.54 and 046, were inaccurate by an amount equal to ±0.06. As theconventional 95% margin of error, this value (0.06) is 1.96 times

$\sqrt{0.5\frac{1 - 0.5}{285}}.$Use of the standard error of measurement would produce a 95% confidenceinterval of the same size, to two decimal places. Substituting the valueof 0.04 for S_(P) and the {circumflex over (ρ)}_(πP) ² value of 0.42 forρ_(πP) ² yields σ_(P-π)=0.03, or (multiplying 0.03 by 1.96) a 95% errormargin of 0.06.

Confidence intervals determined from the standard error of measurementare directly comparable to conventionally determined confidenceintervals because both are based on the assumption of a fixed π and avariable P. The standard error of estimate, applicable in the regressioncase of a fixed P and a variable π, has the same formula as the standarderror of measurement with the exception that σ_(π) replaces S_(P):σ_({circumflex over (π)}−π)=σ_(π)√{square root over (1−ρ_(πP) ²)}. Sinceρ_(πP) ²=σ_(π) ²/s_(P) ², the standard error of estimate will, exceptwhen ρ_(πP) ²=1, be smaller than the standard error of measurement by afactor of ρ_(πP). Conceptually, the standard error of estimate should besmaller than the standard error or measurement because the difference(P−π) contains a varying component representing bias that is absent inthe difference ({circumflex over (π)}−π). Credibility intervals for avariable π will therefore generally be smaller than correspondingconfidence intervals for a variable P. When K=2, the estimate ofσ_({circumflex over (π)}−π) corresponding to σ_(p-π) is equal to theestimate of σ_(p-π) multiplied by {circumflex over (ρ)}_(πP).

In the Carnation example, with {circumflex over (ρ)}_(πP) ²=0.42, thestandard error of estimate is √{square root over (0.42)} times 0.03 (thestandard error of measurement), or 0.019, so that the 95% margin oferror (1.96 times 0.019) is ±0.04. This (rounded from 0.037) isconsiderably smaller than the conventional error margin of ±0.06. Thesample of 285 would, in fact, have to be 417 larger (a total of 702respondents) to achieve the same ±0.04 margin of error conventionally.Since both the confidence and the credibility intervals overlap thechance proportion of 0.50, the data do not support the claim thattasters can tell cream from Coffee-mate.

For the ±credibility interval to be comparable to the conventional ±0.06confidence interval, it must also contain 95% of the area under itsfrequency curve. The next section will investigate the extent to whichthis is the case not only here but also more generally.

Question—do ±1.96 Standard Errors Constitute 95% Credibility Intervalsin π-On-P Estimation?

The answer, generally, is yes, as this section demonstrates.

Corresponding to the assumption of a binomial distribution for sampleproportions is the assumption of a beta distribution for populationproportions. One distribution is the conjugate of the other. Dirichletdistributions are correspondingly conjugates of multinomialdistributions. Such assumptions of conjugate distributions are common inBayesian analysis (e.g., Good, 1965, Chapter 3). Under thebeta-distribution assumption, not only does the ±0.04 credibilityinterval of the Carnation example contain 95% of the area under itsfrequency curve but also, as FIG. 6 shows,±1.96σ_({circumflex over (π)}-π) credibility intervals ranging from±0.01 to ±0.1 of point estimates ({circumflex over (π)}) between 0.05and 0.50 will generally contain 95% of possible π values. (No subscriptfor π is necessary here because a beta distribution involves only twoproportions, π and 1−π.) The coverage shown in FIG. 6 is based oncalculations, not Monte Carlo sampling. FIG. 6 shows confidence-intervalproportions for values of {circumflex over (π)} not only between 0.05and 0.50 but also, though indirectly, between 0.50 and 0.95. Becausebeta distributions having mean values between 0.50 and 0.95 are mirrorimages of beta distributions having mean values between 0.05 and 0.50,the credibility-interval proportion of {circumflex over (π)} is equal tothe credibility-interval proportion for 1−{circumflex over (π)},provided that both distributions have equal standard deviations(σ_({circumflex over (π)}−π)).

FIG. 6 shows credibility-interval proportions as a function ofbeta-distribution means and standard deviations because these are theparameters involved in the determination of credibility intervals. Betadistributions, however, are functions of two parameters, a and b,related to beta-distribution means ({circumflex over (π)}) and standarddeviations (σ_({circumflex over (π)}−π)), as follows:

$\begin{matrix}{{a = {\hat{\pi}\left( {a + b} \right)}}{and}} & (20) \\{{b = {\left( {1 - \hat{\pi}} \right)\left( {a + b} \right)}}{where}} & (21) \\{{a + b} = {\frac{\hat{\pi}\left( {1 - \hat{\pi}} \right)}{\sigma_{\hat{\pi} - \pi}^{2}} - 1}} & (22)\end{matrix}$For the Carnation data, a+b=0.52(1−0.52)/(0.019)²−1, or 690, so thata=0.52(690), or 359, and b=(1−0.52)690, or 331, and for these values ofa and b the interval between 0.52−1.96(0.019) and 0.52+1.96(0.019)contains 95% of the area under the beta-distribution frequency curve.FIG. 6 shows this result in the row for 0.50 (close to 1-0.52) and thecolumn for 0.020 (close to 0.019).

The two standard errors, the standard error of measurement and thestandard error of estimate, differ not only in the lengths of theconfidence or credibility intervals that they produce but also in oneother important respect: While the standard error of measurement,computed from S_(p) and {circumflex over (ρ)}_(πP) ² (itself a functionof S_(p)), is subject to sampling variation due to the possiblevariation of P for each option over samples, the standard error ofestimate does not change because under the π-on-P assumption governingits use each P remains constant while only π can change for each option.In both these respects, the standard error of estimate is superior tothe standard error of measurement for use in the determination ofcredibility or confidence intervals for population proportions.

The results shown in FIG. 6 are applicable to Dirichlet as well betadistributions because a beta distribution describes each Dirichletproportion if all the other proportions are aggregated as itscomplement.

Step 4B—Interval Estimation: Case of Two or More Options

The standard error of estimate provides the basis for a commoncredibility interval, applicable particularly for proportions when K=2.The assumption of a Dirichlet prior distribution for the populationproportions makes possible the determination of a different interval foreach proportion when K≥3. (When K=2, the two intervals are the same.) Ifτ designates the total of the parameters of a Dirichlet distribution,then the posterior variance of π_(k) for option k is

$\begin{matrix}{{{Var}\left( \pi_{k} \middle| P_{k} \right)} = \frac{{\hat{\pi}}_{k}\left( {1 - {\hat{\pi}}_{k}} \right)}{\tau + n + 1}} & (23)\end{matrix}$where P_(k) is the observed proportion for option k. Use of thisequation, with {circumflex over (π)}_(k) from Equation 19 estimating theexpected value of π_(k), requires knowledge of τ.

Since

$\begin{matrix}{\mspace{20mu}{{E\left( \pi_{k} \middle| P_{k} \right)} = \frac{{\hat{\tau\;\pi}}_{k} + {nP}_{k}}{\tau + n}}} & \mspace{14mu}\end{matrix}$for a Dirichlet distribution, the Direchlet shrinkage coefficientcorresponding to {circumflex over (ρ)}_(πP) ² here is n/(τ+n), and sofor the π-on-P procedure

$\begin{matrix}{\tau_{\pi|P} = \frac{n\left( {1 - {\overset{\Cap}{\rho}}_{\pi\; P}^{2}} \right)}{{\overset{\Cap}{\rho}}_{\pi\; P}^{2}}} & (24)\end{matrix}$or, from Equation 15,

$\begin{matrix}{\tau_{\pi|P} = \frac{K\left( {1 - {\sum\limits_{k = 1}^{K}P_{k}^{2}}} \right)}{{K{\sum\limits_{k = 1}^{K}P_{k}^{2}}} - 1 - \frac{K - 1}{n}}} & (25)\end{matrix}$According to Fienberg and Holland (1973), the minimax value of τ is√{square root over (n)} and the maximum-likelihood estimator of τ is

$\begin{matrix}{\tau_{\pi|P} = \frac{K\left( {1 - {\sum\limits_{k = 1}^{K}P_{k}^{2}}} \right)}{{K{\sum\limits_{k = 1}^{K}P_{k}^{2}}} - 1}} & (26)\end{matrix}$

Using the values of T computed from Equations 25 and 26 as well as{circumflex over (π)} in Equation 23 with data from the three-textbookexample presented earlier produced the 95% credibility or confidenceintervals in FIG. 7, which includes corresponding traditional,frequentist results.

FIG. 7 shows separate error margins for the three different textbookgroups, as well as their different population-proportion estimates underthe four different procedures, corresponding in the case of thenon-frequentist procedures to their different values of τ: 21, 232, and487, respectively, for minimax, Fienberg and Holland, and π-on-P. Theπ-on-P procedure produced the narrowest margins of error. This result isnot surprising since, as Equations 25 and 26 make clear, the τ_(π|P) forthe π-on-P procedure is greater than the τ_(π|P) for themaximum-likelihood procedure, the difference diminishing as n getslarge. According to Equation 24, the τ for the minimax procedure(√{square root over (n)}) will also be smaller than the τ_(π|P) for theπ-on-P procedure unless

$\mspace{20mu}{{{{\hat{\rho}}_{\pi\; P}^{2}/\left( {1 - {\hat{\rho}}_{\pi\; P}^{2}} \right)} \geq \sqrt{n}},}$which is not the case here. Since the standard error of the threenon-frequentist procedures ranged between 0.01 and 0.02, the credibilityintervals, as FIG. 6 shows, all have approximately 95% coverage.

Depending on which of the four estimation procedures they use,investigators looking at the study's result might come to entirelydifferent conclusions. Since 0.33 bordered or lay outside the confidenceor credibility interval of the older textbook in all but the π-on-Pprocedure, investigators using the three other procedures might concludethat the older textbook was less effective than the two new ones. Aninvestigator using the π-on-P procedure, however, would not reach thatconclusion. All three credibility intervals produced by that procedure,despite being generally narrower than the other by 0.02, contain thechance proportion 0.33. The conclusion following from this result isthat further study is necessary before selecting a textbook for generaluse.

Comparison of Bayesian and Frequentist Procedures in the Two-Option Case

FIGS. 3 and 4 provide a broader view of the frequentist and Bayesian(π-on-P) regression procedures. Limited to K=2, these figures show, fora range of sample sizes and P values, Bayesian margins of error aspercents of frequentist margins of error (FIG. 3) and actual Bayesianmargins of error (FIG. 4). The three high curves in each figurerepresent high {circumflex over (ρ)}_(πP) ² values, approaching one forn=500, while the bottom curve (for P=0.54) represents a comparably low{circumflex over (ρ)}_(πP) ² value. When {circumflex over (ρ)}_(πP) ² ishigh, Bayesian and frequentist margins of error are very nearly equal,as are corresponding point estimates; when {circumflex over (ρ)}_(πP) ²is low, Bayesian margins of error are low and point estimates are closeto the mean, relative to their frequentist counterparts. For P valuesthat are very close to the mean, {circumflex over (ρ)}_(πP) ² can be solow that Bayesian point estimates are for all practical purposes equalto the mean, with margins of error effectively equal to zero. This isthe case for P=0.52 when n≤500.

Is π-On-P an Empirical or a Purely Bayesian Procedure?

In regressing the observed proportion P toward the mean, 1/K, thesquared correlation {circumflex over (ρ)}_(πP) ² resembles the shrinkagecoefficient w in Fienberg and Holland (1973) or 1−B in Efron and Morris(1973) and Morris (1983). Because the development using w or 1−Binvolves empirical Bayesian estimation, this resemblance suggests thatπ-on-P regression may also be empirical Bayesian. This is not the case,however.

The π-on-P procedure is a regression, not an empirical Bayesian,procedure. The difference is important. While estimates in both theπ-on-P and the Fienberg and Holland (1973) procedures are expectedvalues of π given P, both π and P may vary in empirical Bayesianestimation while only π may vary in estimation by π-on-P regression. IfP as well as π were to vary in π-on-P regression, then the credibilityintervals computed from the standard error of estimate would be toosmall, as the coverage proportions in FIG. 6 would be too large. Themore apt comparison is with pure Bayesian estimation because in boththis and π-on-P regression P is fixed while only π may vary. Because theregressed-on variables are fixed in regression estimation, the coverageproportions in Table 1 and their corresponding credibility intervals areaccurate.

Shrinkage Coefficients as Slope Coefficients in Regression

Shrinkage coefficients may be interpretable as slope coefficients inregression. Stigler (1990) made this observation in relation to the workoriginated by James and Stein (1961), involving means. The squaredcorrelation ρ_(πP) ² is in fact the single-sample binary (0-1) datacounterpart to the multi-sample shrinkage coefficient 1−B cited byMorris (1983) under the empirical Bayes assumption of normaldistributions for both sample and population means. If for m samples,with m>2, σ _(X−μ) ² and σ_(π) ² are the respective variances of thesedistributions, then according to Morris,

$\mspace{20mu}{{{1 - B} = {\sigma_{\mu}^{2}/\left( {\sigma_{\mu}^{2} + \sigma_{\overset{\_}{X} - \mu}^{2}} \right)}},}$which is the square of the correlation between μ and X. In view ofEquations 23 and 24, the shrinkage coefficient w in the Fienberg andHolland (1973) procedure is a consistent estimator of the squaredcorrelation between n and P, the shrinkage coefficient developed here({circumflex over (ρ)}_(πP) ²) being an unbiased estimator of it.

In the James-Stein case, σ _(X−μ) ²=1, so that the risk function of thetraditional estimator X is equal to one for each value of μ. If μ has anormal distribution, then the posterior variance of μ given X is equalto 1−B and, as Efron and Morris (1973) observed, the risk function ofthe estimator (1−B)X, which is the posterior mean of μ given X assumingμ to have a mean of zero, is smaller than the risk function of X by anamount equal to B. The estimator (1−B)X is the James-Stein estimator ifB is replaced by

${\left( {m - 2} \right){\sum\limits_{i = 1}^{m}{\overset{\_}{X}}_{i}^{2}}},$whose expected value is equal to B because, with σ _(X−μ) ²=1 and 1/Bthe variance of X,

$\sum\limits_{i = 1}^{m}{\overset{\_}{X}}_{i}^{2}$has a χ² _(m) distribution with negative first moment equal to 1/(m−2).The James-Stein shrinkage coefficient

$1 - {\left( {m - 2} \right){\sum\limits_{i = 1}^{m}{\overset{\_}{X}}_{i}^{2}}}$is thus interpretable as an unbiased estimator of the square of thecorrelation between μ and X, or as an unbiased estimator of the slopecoefficient in the regression of μ on X, the intercept being equal tozero.

Since {circumflex over (ρ)}_(πP) ² is also an unbiased estimator, itcorresponds in single-sample proportions estimation to the James-Steinshrinkage coefficient in multi-sample means estimation.

Bayesian Versus Conventional Margins of Error in the Estimation ofPopulation Proportions

The use of Bayesian (π-on-P) estimation of a population proportionrequires an amended definition of margin of error. In the conventionalor classical estimation, the margin of error depends only on samplesize. In Bayesian estimation, the margin of error varies not only withsample size but also with the estimated population proportion obtainedfrom the observed sample proportion. The margin of error that ismeaningful in Bayesian estimation is the difference between 0.50 and theestimated population proportion. This margin of error is called thecritical margin of error. When the estimated population proportion is0.53, for example, the critical margin of error is 0.03 (0.53−0.50). If0.53 is the Bayesian estimate of the population proportion in a samplelarge enough to produce a margin of error equal to 0.03 for aBayesian-estimated population proportion of 0.53, then the conclusionfrom this result is that the population proportion is equal to 0.53 plusor minus 0.03 or, in other words, that the population proportion ismarginally larger than 0.50. If for the same sample size the Bayesianpopulation proportion estimate is larger than 0.53 (or smaller than0.47), then that estimate plus or minus its error margin will exclude0.50, as illustrated by the accompanying graph of a sample of size 300(FIG. 8).

For conventional and Bayesian estimation of population proportions, FIG.9 shows a table that compares samples sizes required to produce commonlyused error margins. In the case of Bayesian estimation, the errormargins are critical margins of error. For every error margin, Bayesianestimation requires a smaller sample than conventional estimation.

So far, this disclosure has shown that the regression of population onsample proportions constitutes Bayesian estimation of populationproportions, while demonstrating how to perform this estimation onexisting data. To apply this procedure to the conduct of a survey, whatis needed is a method to determine the sample size n required to achievea desired margin of error. In conventional or classical estimation, themethod involves the use of a simple formula: n=0.96/m², where mdesignates the desired margin of error. For a margin of error equal to0.03, for example, n=0.96/(0.03), or 1,067. This simple determination ofsample size is possible because n is a function only of m inconventional estimation. Unfortunately, in Bayesian estimation thesample size is a function of two variables, the desired margin of errorand the yet-to-be-observed sample proportion, which is itself a functionof ρ_(πP) ² and n, as shown in the following equation for ρ_(πP) ² withK=2:

$\begin{matrix}{\mspace{20mu}{\rho_{\pi\; P}^{2} = {2\left( {\sqrt{\left( {\left( {n - 1} \right)m^{2}} \right)^{2} + {n\; m^{2}}} - {\left( {n - 1} \right)m^{2}}} \right)}}} & (27)\end{matrix}$

In conventional estimation, the margin of error is the distance betweentwo boundaries that contain between them 95 percent of outcomes thatcould occur by chance. That is necessary because in this estimation thechance outcomes could occur with equal probability on either side of thenull population value (equal to 0.50 when estimating probabilities withK=2). The occurrence of an observation outside the error marginnullifies that population value (meaning, when K=2, it is not equal to0.50). The same necessity does not exist in Bayesian estimation,however, because in this case the observed value is considered fixed,meaning it cannot be on the other side of the null value, and the errormargin needs only a boundary between it and the null value, the distancebetween the estimated value and the boundary being the margin of error(0.03, for example). Nullification occurs when the null value (0.50 whenK=2) is outside this error margin. A one-sided error margin has adistinct advantage over a two-sided one, which divides the 5 percentnullification values into two equal 2.5 percent parts, one on eitherside. In the case of a one-sided error margin, all the 5 percentnullification values are on one side, the side where the null value is.A sample that has 5 percent of nullification values on a single side ofan error-margin boundary must be smaller than one that has 2.5 percentthere. That means a smaller sample will produce any desired margin oferror in the one-sided than in the two-sided case. This sample-sizeadvantage is in addition to the one that Bayesian estimation has even inthe case of two-sided error margins. FIG. 9 provides an illustration forselected margins of error.

Substituting a constant value (like 0.03) for m and successive integervalues of n in this equation, while at each simulation using theresulting value of ρ_(πP) ² to determine the actual margin of error,until reaching the value of n that yields an actual error margin that isequal to the desired one (m) leads to the determination of the requiredsample size. This process requires the use of a computer.

In other embodiments of the regression of a true measurement on anobserved measurement to be described below, the true measurement may bethe probability of a correct response to an item by a particular personand the observed measurement may be an actual response by that person tothat item. The probability of a correct response regressed on anobserved measurement may be implemented in a delta single-parameterlogistic modeling technique. Delta single-parameter logistic modelingincludes receiving a response for each item and each test taker, and atotal score for each test-taker. The correlation between the itemresponse and the total test score or other ability measure over theplurality of test-takers is determined. A Bayesian-estimated probabilityof a correct response for each test-taker to each item is determined.The logit of this probability estimate is determined. The difficulty ofan item is estimated as a function of the average logit over the numberof test-takers. A test-taker's ability is likewise estimated as afunction of the average logit and the average difficulty over the numberof test items.

Referring now to FIGS. 10A and 10B, a delta single-parameter logisticmodeling method, in accordance with a number of embodiments of thepresent technology, is shown. It is appreciated that method may beimplemented in hardware, software, firmware and/or a combinationthereof. The method may begin with receiving correct or incorrect itemresponses (x) for each test taker and each test item, and a total testscore (X) for each test-taker, at 110. In one implementation, the testmay be a multi-item test taken by a plurality of test takers. In oneimplementation, the value of the response (x) may be 1 if correct and 0if incorrect.

At 115, a correlation (ρ) between the item response (x) and the totaltest score (X) over the plurality of test takers is determined. Thecorrelation (ρ) incorporates item discrimination, which is the extent towhich an item measures what it is supposed to measure (what the test asa whole measures). At 120, an initial Bayesian estimate of theprobability (P) of a correct response for each test-taker to each itemis determined as the weighted average of x and the proportion oftest-takers who answer the item correctly, wherein the weights are ρ²for x and one minus ρ² for the proportion of test-takers who answer theitem correctly.

At 125, a logit (λ) of the probability (P) is determined, wherein thelogit of a probability is the natural logarithm of the ratio of theprobability to it complement (i.e., one minus the probability). At 130,a difficulty of an item is estimated as minus the average logit (λ) overthe number of test-takers. At 135, a given test-taker's ability (θ) isestimated as the sum of the average difficulty (b) and the average logit(λ) over the number of test items. At 155, the difficulty of the itemand the test-taker's ability are output. In one implementation, theresults may be output by storing on a computing device readable medium(e.g., computer memory), displaying on a monitor (e.g., computerscreen), and/or the like.

At 140, the processes of 115-135 may optionally be iteratively repeated,replacing the correlation between item response and total test scorewith the correlation between item response (x) and test-taker ability(θ), one or more times or until a change in the given test-taker'sability (θ) is less than a predetermined amount. In one implementation,the processes of 115-135 may be iteratively repeated until no estimateof given test-taker's ability (θ) differs from its estimate on thepreceding cycle by more than 0.004. In such an implementation, a giventest-taker's ability (θ) estimates are used until they settle intostable values, when the iterative process ends.

At 145, new difficulty (b) estimates for use in item banks and test-formequating may optionally be determined using a simple Rasch model. At150, the process is to retain these Rasch b estimates while adding theiraverage to each θ estimate obtained in processes 115-140 and subtractingfrom it the average b estimate obtained in these processes (115-140).

The processes of 115-135 are referred to herein as the deltasingle-parameter logistic model. The processes of 115-140 implement thedelta single-parameter logistic model providing an improved test-taker'sability (θ). The processes of 115-150, which combine the deltasingle-parameter logistic model for the test taker's ability (θ) and theRasch model for the item difficulty (b), is referred to herein as thehybrid model.

Embodiments of the present technology will be further elucidated in thefollowing description, which will begin with data adjustment, movingfrom the test level to the item level via the point-biserial correlationr_(xX) between item response (x) and test score (X). In classical testtheory, this correlation, with the difference between the mean testscore of examinees who get the item right and the mean test score ofexaminees who get the item wrong in the numerator (McNemar, 1962, p.192), not only measures but in fact captures the precise meaning of itemdiscrimination. The section following the next will describe parameterestimation in the fit of the Rasch model to item-response data adjustedby r_(xX). (Throughout this description, lower-case x will refer to anitem and upper-case X to a test.)

Logits: Observed Scores Vs. True-Score Estimates

The logit of a proportion or probability P is the natural logarithm ofthe ratio of P to (1−P). If P_(iq) is the probability of a correctresponse to item i by examinee q, then for a two-parameter logisticmodel the logit of P_(iq) is equal to a_(i)(θ_(q)−b_(i)). The nextsection will need an empirical counterpart of P_(iq) that has acomputable logit. In the Gulliksen version of classical test theory, theobserved response x_(iq), equal to zero or one, qualifies as such acounterpart, but x_(iq) has no computable logit. The Kelley version ofclassical test theory suggests a solution to this problem: a weightedaverage of an observed score and the observed score mean for a singleitem,

$\begin{matrix}{p_{iq} = {{r_{xX}^{2}x_{iq}} + {\left( {1 - r_{xX}^{2}} \right){\overset{\_}{x}}_{i}}}} & (27)\end{matrix}$where x _(i) is the item difficulty, equal to the mean of x_(iq) over q,and r_(xX), as noted earlier, is a measure of item discrimination.Unless r_(xX) is equal to one, which is not possible for apoint-biserial correlation, or xi is equal to zero or one, P_(iq) willalways have a computable logit.

The use of P_(iq) rather than x_(iq) as the empirical counterpart ofP_(iq) makes rational as well as mathematical sense, especially formultiple-choice items. On a zero-to-one scale, the value of one isarguably too large a measure for a correct item response in this case.Not only may the knowledge or skill measured by an item be at leastpartially irrelevant but also a correct response does not necessarilyreflect that knowledge or skill. Guessing may play a role. The effect ofguessing, however, also depends on the item's discrimination. Anexaminee who gets the item right is more likely to have a high testscore and thus be more generally knowledgeable on the subject tested ifthe item is highly discriminating than otherwise. This difference isproperly reflected in the value of p_(iq). If x_(iq)=1 for each of twoitems that vary in discrimination, the value of p_(iq) will be closer toone for the more discriminating than for the less discriminating item. Alike argument applies in the case of incorrect answers to the two items.A correct answer has a greater positive effect and an incorrect answer agreater negative effect on p_(iq) if made to a more discriminating thanto a less discriminating item in contrast to the uniformly equalpositive and negative effects on x_(iq). For a multiple-choice item andarguably for any item scored on a zero-to-one scale, p_(iq) willgenerally be a more precise reflection of relevant item knowledge thanx_(iq).

Fitting the Model to Data

Transformation of the single-parameter logistic model of Equation (1) toa logit form permits the accommodation of item discrimination anddifficulty in model fit. With the addition of the indices i and q,Equation (1) has the logit form

$\begin{matrix}{{\ln\left( \frac{P\left( {\theta_{q};b_{i}} \right)}{1 - {p\left( {\theta_{q};b_{i}} \right)}} \right)} = {\theta_{q} - b_{i}}} & (28)\end{matrix}$the left side corresponding empirically to ln (P_(iq)/[1−P_(iq)]), thelogit of p_(iq). If λ_(iq)=ln (P_(iq)/[1−P_(iq)]), then empirically thelogit form of the single-parameter logistic model is a regressionequation like X=T+E of classical test theory, ∈_(iq) denoting error:

$\begin{matrix}{\lambda_{iq} = {\left( {\theta_{q} - b_{i}} \right) + ɛ_{iq}}} & (29)\end{matrix}$so that, as ordinary least-squares estimates assuming the average θ tobe equal to zero,

$\begin{matrix}{{{\hat{b}}_{i} = {- \overset{\_}{\lambda_{i}}}}{and}} & (30) \\{{\hat{\theta}}_{q} = {{\overset{\_}{\lambda}}_{q} + \overset{\_}{b}}} & (31)\end{matrix}$λ _(ι) being the mean of λ_(iq) over q, λ _(q) the mean of λ_(iq) overi, and b the mean item difficulty. All computation of item statistics inthis and subsequent sections are over q.

Since {circumflex over (b)}_(i) is a mean and θ_(q) a sum of means,estimates of their sampling variances are, respectively,

$\begin{matrix}{\mspace{70mu}{{{\hat{\sigma}}_{b_{i}}^{2} = \frac{\sum\limits_{q = 1}^{N}\left( {\lambda_{i\; q} - {\overset{\_}{\lambda}}_{i}} \right)^{2}}{N\left( {N - 1} \right)}}\mspace{20mu}{and}}} & (32) \\{= {\frac{\sum\limits_{i = 1}^{N}\left( {\lambda_{i\; q} - {\overset{\_}{\lambda}}_{q}} \right)^{2}}{n\left( {n - 1} \right)} + \frac{\sum\limits_{i = 1}^{n}{\hat{\sigma}}_{b_{i}}^{2}}{n^{2}}}} & (33)\end{matrix}$

Despite their apparent equivalence, differing only by a scaleconversion, ϑ_(q) and (ϑ_(q)−b_(i)) cannot both be true scores, like T.Just as a true score in classical test theory must be both test- andexaminee-dependent, so a true score for an item in modern test theorymust be both item- and examinee-dependent. This is the case for(ϑ_(q)−b_(i)), but not for ϑ_(q), which is only examinee-dependent,though both θ_(q) and (ϑ_(q)−b_(i)) have equal standard deviations andcorrelations with other variables.

For a single sample, in fact, the correspondence between T andϑ_(q)−b_(i) is striking if, following common practice, the mean of θ isassumed to be zero. In this case, just as the mean of T (T), being equalto the proportion of examinees (X) who get the item right, is themeasure of item difficulty in classical test theory, so the mean ofϑ_(q)−b_(i) over examinees is equal to minus the measure of itemdifficulty (b_(i)) in modern test theory. While X decreases, minus themean of ϑ_(q)−b_(i) over examinees increases with increasing itemdifficulty.

The accommodation of varying item discrimination by using thetrue-response estimates p_(iq) to estimate b_(i) and θ_(q) occurswithout the use of the parameter α_(i). The resulting single-parameterlogistic model retains the parameter-separation but not theunweighted-scores property of the Rasch model, a price paid for by theaccommodation of varying item discrimination. As weighted scores, thetrue-response estimates p_(iq) reflect varying item discrimination incontrast to the observed responses x_(iq) which do not. The combinationof a weighted average like the Kelley estimation model of classical testtheory with the Rasch model of modern test theory extends the usefulnessof the single-parameter logistic model empirically to tests consistingof items that vary substantially in discrimination.

The Parameter α and Item-Test Correlation

In classical test theory, as noted earlier, the correlation between theresponse x (0 or 1) to an item and total test score X provides a measureof the discrimination of the item. According to Lord and Novick (1968,p. 378, Equation 16.10.7), in the case of the two-parameter normal-ogivemodel with θ assumed to have a standard normal distribution (μ_(θ)=0 andσ_(θ)=1), the slope parameter α has the following relationship to acorrelation similar to r_(xX): α=r_(xθ)/√{square root over (1−r_(xθ)²)}, where r_(xθ) is a biserial correlation between θ and x equal toeither zero or one depending on the value of a latent item-knowledgevariable (y) having a standard normal distribution (μ_(y)=0 andσ_(y)=1). The numerator in the formula for this correlation, like thenumerator in the corresponding formula for r_(xX), involves thedifference between the mean θ of examinees who get the item right andthe mean θ of examinees who get the item wrong (McNemar, 1962, p. 189).Since r_(xθ)/√{square root over (1−r_(xθ) ²)} is a direct function ofr_(xθ), α in the two-parameter normal-ogive model is a measure of itemdiscrimination in the tradition of classical test theory.

In this tradition, Birnbaum (1968, pp. 402-403) in fact justified theneed to measure discrimination in item response models by citingstandard deviations of actual item-test correlations (r_(xX)) that weretoo large to occur by chance. This illustration is of interest here fortwo reasons: Birnbaum recognized (a) the near-equivalence of X and θ and(b) the role of dichotomous item response measures (x) in themeasurement of item discrimination (“ . . . item-test biserialsapproximate item-ability biserials . . . ”). The remainder of thissection examines the relationship between item discrimination and theparameter a in Birnbaum's two-parameter logistic model.

Like the normal-ogive model, the two-parameter logistic model must maketwo assumptions in order to have α_(i) as a slope parameter of(θ_(q)−b_(i)). A straightforward way to expose the two assumptions inthis case is to change the scale of 0 by multiplying(θ_(q)−b_(i))+ε_(iq) through by the value of α_(i) (different from aregression coefficient) that will make the standard deviation of theproduct of α_(i) and ε_(iq) equal to one (S_(α) _(i) _(ε) _(iq) =1)while assuming, as is traditionally the case, that S_(θ)=1. Just asX=T+E in classical test theory implies that S_(T)=r_(XT)S_(X), so inmodern test theory λ_(iq)=(θ_(q)−b_(i))+ε_(iq) implies thatS_(θ)=r_(λθ)S_(λ). Since S_(α) _(i) _(ε) _(iq) =1 andS_(θ)=r_(λθ)S_(λ)=1,

$\begin{matrix}{{\alpha_{i}ɛ_{iq}} = {\frac{ɛ_{iq}}{s_{ɛ_{iq}}} = {{\left( \frac{1}{S_{\lambda}\sqrt{1 - r_{\lambda\;\theta^{2}}}} \right)ɛ_{iq}} = {{\left( \frac{r_{\lambda\theta}S_{\lambda}}{S_{\lambda}\sqrt{1 - r_{\lambda\;\theta}^{2}}} \right)ɛ_{i\; q}} = {\left( \frac{r_{\lambda\;\theta}}{\sqrt{1 - r_{\lambda\;\theta}^{2}}} \right)ɛ_{iq}}}}}} & (34)\end{matrix}$which shows that the value of α_(i) corresponds to the value presentedby Lord and Novick (1968, p. 378) and by Lord (1980, pp. 31-32) in thenormal-ogive case: r_(λθ)/√{square root over (1−r_(λθ) ²)}, λ like xbeing a two-valued variable measuring the response to an item. (In fact,r_(λθ)=r_(λθ).) In Equation (12), the second value

$\left( \frac{ɛ_{iq}}{s_{ɛ_{iq}}} \right)$is the standardized value of ε_(iq), whose standard deviation is equalto one. The two assumptions are S_(α) _(i) _(ε) _(iq) =1 and S_(e)=1.Iteration from r_(xX) to r_(xθ)

The near-equivalence of X and θ cited by Birnbaum (1968) suggests theuse of iteration to estimate θ_(q) and b_(i) by way of λ_(iq). The firststep in the iteration is to use r_(xX) to estimate λ_(iq), as describedearlier. Subsequent steps use r_(xθ), the θ values being obtained fromthe preceding step. Iteration continues till the estimates of both θ andb stabilize. Since r_(xθ)=r_(x(θ−b)), the result of this procedurecorresponds precisely at the item level to Kelley's at the test level,(θ−b) and T both being true scores, as indicated earlier. Since theparameter a is a function of r_(xθ) or r_(λθ) (the two being equal), theuse of r_(λθ) to measure item discrimination is tantamount to the use ofα to do so in the two-parameter normal-ogive and logistic models. Testanalysts who wish to follow the Verhelst and Glas (1995) approach to theaccommodation of varying item discrimination in a single-parameterlogistic model can also use r_(λθ) in α=r_(xX)/√{square root over(1−r_(xX) ²)} or r_(xθ) in α=r_(xθ)/√{square root over (1−r_(xθ) ²)} tohelp determine the “constant” value of α in their model. The second ofthese is the equation used in the following numerical example.

Numerical Example

This section describes the use of simulated data to compare thesingle-parameter model in its traditional Rasch form with the formdescribed here, as well as a number of variations of each. The dataconsisted of individual item responses (0's and 1's) on 10-, 20-, and30-item tests, each administered to 1,000 examinees. The θ values wererandomly selected from a standard normal distribution. The b values forthe 10-item test were −1.5, −0.75, 0, 0.75, and 1.5, each repeated once.These ten values were duplicated in the 20-item test and triplicated inthe 30-item test. Items created to have each b value were also createdto have either of two values of r_(xθ). One is the maximum possiblevalue for its difficulty, point-biserial correlations having maximumvalues less than one, and the other is 0.144, chosen to make the meanr_(xθ) for a test equal to 0.400. Table 1 shows these r_(xθ) values inthe third-to-last row.

Data creation. Using the 0 values and item specifications justdescribed, the following regression equation yielded the probability ofa correct response, P_(iq), for each item and each examinee:P_(iq)=√{square root over (P _(i)(1−P _(i)))}r_(xθ)θ_(q)+P _(i), where P_(i) is the mean over examinees of 1/(1+e^(−(θ) ^(q) ^(−b) ^(i) ⁾). Thisequation for P_(iq) accommodates both item difficulty (P _(i)) and itemdiscrimination (r_(xθ)) in the determination of the probability of acorrect response. Comparison of P_(iq) with R_(iq), a random numberuniformly distributed between 0 and 1, resulted in the simulatedresponse: 1 if R_(iq)≤P_(iq), 0 otherwise. This procedure determined theresponse of each examinee to each item on each of the three tests.

Comparison of models. To facilitate comparison with other models, themodel described here will be called the delta model because it involvesan increment to the single-parameter logistic model to account for itemdiscrimination. In addition to the Rasch model, these other models arethe Rasch K model, a single-parameter model described by Verhelst andGlas (1995) that has a “constant” (unfitted) discrimination parameter(α) that may differ in value from item to item, the biserial delta modelthat uses biserial instead of point-biserial correlations for r_(xθ) inthe delta model, and a hybrid model involving Rasch model estimates of band delta model estimates of θ. The focus of the comparisons will be onthe correlations of b and θ estimates with their true values.

Estimation procedures. The estimation procedures differed for the Raschand the delta models. Estimation for the Rasch and Rasch K models usedthe maximum-likelihood procedure described by Wright and Panchapakesan(1969), with Newton-Raphson iteration. Involving least squares,estimation for the delta and biserial delta models used Equations (8)and (9) together with the iteration procedure described in the precedingsection. Iteration continued till the difference between successiveestimates was equal to zero, to two decimal places. The “constant”discrimination parameter a used in the Rasch K model for each item wasequal to r_(xθ)/√{square root over (1−r_(xθ) ²)}, the θ values being theones estimated separately for each test with the delta model. Both theRasch and the delta estimation procedures involved joint (b and θ)estimation without any distributional assumptions.

Estimation for the hybrid model used the Rasch procedure for b and, withthe b values fixed at their Rasch estimates, the delta procedure for θ.

Results. The five models lined up differently for the estimation of band the estimation of θ. One tied at the top for b but not for θ, onetied at the top for θ but not for b, and one tied at the top for both.Two did not fare well for either b or θ. The b comparisons involved onlythe items in the 10-item test, the items in the other tests beingspecification replicates of these items. The θ comparisons involved allthree tests.

Referring now to the table in FIG. 11, the b results are shown. The toppart of the table displays the difficulty and the bottom part thediscrimination data for the ten items. The fourth row, labeled “True” inthe left column, contains the values of b and the third-from-bottom rowcontains the values of r_(xθ) used to create the item responses. Theremaining rows contain data obtained by parameter estimation. Thenumerically labeled columns in the top part of the table display the bestimates ({circumflex over (b)}) for the ten items obtained from theuse of each of the five models, as well as the mean of these estimatesand the correlation, r_({circumflex over (b)}b), between them and theirtrue values. The r_({circumflex over (b)}b) correlations aresubstantially lower for the biserial delta, Rasch K, and delta modelsthan for the Rasch and hybrid models, being close to one for the lattertwo. The mean b estimates are also substantially farther from the truemean for the delta, biserial delta, and Rasch K models than for theRasch and hybrid models, whose common mean (−0.01) is very nearly equalto the true mean (0.00). The row labeled {circumflex over (α)} containsthe α estimates for the ten items, along with their mean. The bottom rowcontains values of the t-Fit statistic described by Wright and Masters(1982, Ch. 5) and Masters (1988). These statistics are interpretablemore or less like t statistics, highly negative values expected foritems of high discrimination and highly positive values expected foritems of low discrimination. The data bore out these expectations, ascomparison of the t-Fit values with the α estimates in thesecond-to-last row and the r_(xθ) values in the third-to-last row shows.

The line-up of the Rasch and delta models was just the opposite for θestimation, the hybrid model holding the same position in bothcomparisons. Measuring internal validity, the correlations of the θestimates with their true values, which increased with test size, weresubstantially higher in all three tests for the delta and hybrid models(0.807, 0.895, and 0.916) than for the biserial delta model (0.790,0.860, and 0.895) and the Rasch (0.740, 0.848, and 0.883) and Rasch K(0.679, 0.790, and 0.859) models. Referring now to FIG. 12, results forthe delta and hybrid models (top curve) and the Rasch model (bottomcurve) are shown. Of particular interest, seen clearly in this figure,is that the delta and hybrid correlation for the 20-item test was higherthan the Rasch correlation for the 30-item test.

Discussion Although the focus here is on the comparison between theRasch model and the delta model, consideration of the internal-validitystandings of the other models is also informative. In yielding unfittedthough reasonably differing slopes for the item response curves, theRasch K model yielded b and θ estimates that had lower internalvalidities than those of the Rasch model. The biserial delta modelsharpened the r_(xθ) differences among the items but, in so doing,distorted both the b and the θ estimates with a consequent degradationin the internal validities of both. In any case, use of either of thesemodels in practice would be expected to produce no better internalvalidities for either b or θ than their Rasch and delta counterparts.

Because a trade-off exists between test length and item discriminationin their effects on the internal validity of a test (the higher theaverage item discrimination the lower the test length needed to achievea specific internal validity), use of a model that weights itemsdirectly according to their discrimination measures should produce θestimates having higher internal validities than use of a model thatdoes not, all else equal. Estimates of θ should, accordingly, have ahigher internal validity when produced by the delta model than whenproduced by the Rasch model from the same test data. That difference isprecisely what the results here show for each of the three tests. Thedifference is in fact so great that, as noted earlier, use of the deltamodel on the 20-item test produced θ estimates that had a higherinternal validity than use of the Rasch model on the 30-item test.

The superiority of the delta over the Rasch model applies only to theestimation of θ. In the estimation of b, the Rasch model is superior tothe delta model with respect to both the internal validities and themeans of the estimates. This difference is especially important for testdevelopers who maintain item banks and use anchor items in equatingtests.

When one model is better for θ and the other for b estimation and whenthe estimation of both are important, as they both are, which model is atest developer or analyst to use? That question motivated the additionof the hybrid model to the study. The parameter separation that existsin both the Rasch and the delta models suggested that the hybrid modelmight share the advantages of both. The results confirmed thisexpectation. Use of the hybrid model in each test resulted in θestimates that had the internal validity of the delta model and bestimates that had the internal validity, as well as the mean, of theRasch model. In the line-up of all five models, the hybrid model wasclearly the best in both θ and b estimation.

To study the effect of differences in item discrimination on parameterestimation, the data were created in an attempt to maximize thesedifferences while maintaining a reasonable mean value. The t-Fitstatistics, shown in the bottom row of the table in FIG. 11, indicatethat this attempt was successful. Although the dispersion of itemdiscrimination measures in actual data is not likely to be so large asin the data studied here, some dispersion will always occur in practice.The differential effects found in this simulation study maycorrespondingly be smaller in the real world, but they will neverthelessexist there. While retaining the Rasch model's advantages in theestimation of item difficulties, use of the hybrid model will make theexistence of real-world differences in item discrimination less of achallenge than it would be with use of the Rasch model.

Test developers or analysts reading the description in thisspecification may wonder why they should fit a single-parameter model toaccommodate varying item discrimination when the option exists simply touse the two-parameter logistic or normal-ogive model for the samepurpose. One important reason may be this: Single-parameter models donot permit the crossing of item response curves that two-parametermodels do. Another reason, at least as important: Like the delta model,two-parameter models may distort the distribution of b estimates in theprocess of increasing the precision of θ estimation. The first is atheoretical, the second a practical problem. The hybrid model avoidsboth.

Embodiments of the present technology can be used in variousapplications of single-parameter logistic models, which yield theprobability of an outcome as a function of the ability of the performerand the difficulty of the job whenever the purpose of these applicationsis to obtain estimates of performer ability and job difficulty. Use ofembodiments of the present technology will result in improved estimationof performer ability and, combined with the Rasch model, improvedestimation of job difficulty, as well. Examples of other applicationsinclude baseball, involving a player's ability to hit and the difficultyof hitting a pitcher, and spelling bees, involving the spelling abilityof a contestant and the difficulty of a word to be spelled. It is,however, appreciated that embodiments of the present technology are notlimited to these examples.

The foregoing descriptions of specific embodiments of the presenttechnology have been presented for purposes of illustration anddescription. They are not intended to be exhaustive or to limit theinvention to the precise forms disclosed, and obviously manymodifications and variations are possible in light of the aboveteaching. The embodiments were chosen and described in order to bestexplain the principles of the present technology and its practicalapplication, to thereby enable others skilled in the art to best utilizethe present technology and various embodiments with variousmodifications as are suited to the particular use contemplated. It isintended that the scope of the invention be defined by the claimsappended hereto and their equivalents.

What is claimed is:
 1. A method comprising: specifying a desired marginof error for a Bayesian point estimate of an actual populationproportion of individuals choosing a particular one of a plurality ofoptions; determining a sample size needed to achieve the desired marginof error by simulating samples of different sizes, each yielding aBayesian point estimate equal to the desired margin of error plus anumber equal to one divided by the number of the plurality of options,the determined sample size being the one yielding the desired margin oferror; obtaining from a population of individuals each choosing one ofthe plurality of options an independent random sample of the sizedetermined to yield the desired margin of error; and estimating theactual population proportion of individuals choosing the particular oneof the plurality of options from a simple Bayesian regression of theestimated population proportion on the computed proportion ofindividuals in the sample choosing the particular one of the pluralityof options using an unbiased classical, non-Bayesian estimator of asquare of the correlation between the population proportions and thesample proportions over the plurality of options as the regressioncoefficient, wherein said sample size is less than a sample sizerequired to achieve said desired margin of error determined without saidsimulating and said estimating.
 2. The method according to claim 1,wherein the square of the correlation between the population proportionsand sample proportions over the plurality of options is a function of(a) a sum of squares of the sample proportions over the plurality ofoptions computed from the sample of the determined size, (b) the numberof the plurality of options, and (c) the determined sample size.
 3. Themethod according to claim 1, further comprising estimating an actualcredibility interval as a function of a posterior variance of thepopulation proportion of individuals choosing the particular one of theplurality of options.
 4. The method according to claim 3, furthercomprising estimating the actual credibility interval as a function of aBayesian point estimate of the actual population proportion and a totalof parameters of a Dirichlet distribution, which itself is a function ofthe determined sample size and the square of the correlation between thepopulation proportions and the sample proportions over the plurality ofoptions.
 5. The method according to claim 3, further comprisingestimating the actual credibility interval as equal to a Bayesian pointestimate of the actual population proportion of individuals choosing theparticular one of the plurality of options plus or minus 1.96 times asquare root of a posterior variance of the population proportion ofindividuals choosing the particular one of the plurality of options,that substantially includes 95 percent of possible values of thepopulation proportion, while being smaller than a correspondingconfidence interval determined from a conventional estimation of apopulation proportion.
 6. The method according to claim 5, wherein,uniquely applicable to Bayesian estimation, an actual credibilityinterval including the Bayesian point estimate of an actual populationproportion and containing 95 percent of the possible values of theactual population proportion of individuals choosing the particular oneof the plurality of options can have only a single boundary located inthe direction from the point estimate toward the mean populationproportion at a distance from the point estimate equal to 1.6457 times asquare root of a posterior variance of the population proportion ofindividuals choosing the particular one of the plurality of options. 7.The method according to claim 6, wherein the sample size required forany desired margin of error is even smaller than it is for a credibilityinterval having two boundaries at a distance on either side of theBayesian point estimate equal to 1.96 times a square root of a posteriorvariance of a population proportion and containing 95 percent of thepossible values of the population proportion.
 8. The method according toclaim 1, wherein the Bayesian point estimate of a population proportionhas a smaller margin of error for any sample size or requires a smallersample size for any desired margin of error than a conventionallyestimated population proportion.
 9. The method according to claim 1,wherein a mean of the population proportions and a mean of the sampleproportions over the plurality of options are each equal to an inverseof the number of the plurality of options.
 10. The method according toclaim 1, wherein the square of the correlation between the populationproportions and the sample proportions over the plurality of optionsfunctions as a shrinkage coefficient.
 11. The method according to claim1, further comprising estimating an actual margin of error for a pointestimate calculated from the simple Bayesian regression of the estimatedpopulation proportion on the computed sample proportion via a Dirichletdistribution formula if the number of the plurality of options is equalto or greater than two.
 12. The method according to claim 1, furthercomprising estimating an actual margin of error for a point estimatecalculated from the simple Bayesian regression of the estimatedpopulation proportion on the computed sample proportion via a formulafor the standard error of estimate if the number of the plurality ofoptions is equal to two.
 13. A non-transitory computer-readable mediumhaving instructions stored thereon that, responsive to execution by anelectronic system, cause said electronic system to perform operations,the operations comprising: specifying a desired margin of error for aBayesian point estimate of an actual population proportion ofindividuals choosing a particular one of a plurality of options;determining a sample size needed to achieve the desired margin of errorby simulating samples of different sizes, each yielding a Bayesian pointestimate equal to the desired margin of error plus a number equal to onedivided by the number of the plurality of options, the determined samplesize being the one yielding the desired margin of error; obtaining froma population of individuals each choosing one of the plurality ofoptions an independent random sample of the size determined to yield thedesired margin of error; and estimating the actual population proportionof individuals choosing the particular one of the plurality of optionsfrom a simple Bayesian regression of the estimated population proportionon the computed proportion of individuals in the sample choosing theparticular one of the plurality of options using an unbiased classical,non-Bayesian estimator of a square of the correlation between thepopulation proportions and the sample proportions over the plurality ofoptions as the regression coefficient, wherein said sample size is lessthan a sample size required to achieve said desired margin of errordetermined without said simulating and said estimating.
 14. Thenon-transitory computer readable media of claim 13 wherein the square ofthe correlation between the population proportions and sampleproportions over the plurality of options is a function of (a) a sum ofsquares of the sample proportions over the plurality of options computedfrom the sample of the determined size, (b) the number of the pluralityof options, and (c) the determined sample size.
 15. The non-transitorycomputer readable media of claim 13 wherein the operations furthercomprise: estimating an actual credibility interval as a function of aposterior variance of the population proportion of individuals choosingthe particular one of the plurality of options.
 16. The non-transitorycomputer readable media of claim 15 wherein the operations furthercomprise: estimating the actual credibility interval as a function of aBayesian point estimate of the actual population proportion and a totalof parameters of a Dirichlet distribution, which itself is a function ofthe determined sample size and the square of the correlation between thepopulation proportions and the sample proportions over the plurality ofoptions.
 17. The non-transitory computer readable media of claim 15wherein the operations further comprise: estimating the actualcredibility interval as equal to a Bayesian point estimate of the actualpopulation proportion of individuals choosing the particular one of theplurality of options plus or minus 1.96 times a square root of aposterior variance of the population proportion of individuals choosingthe particular one of the plurality of options, that substantiallyincludes 95 percent of possible values of the population proportion,while being smaller than a corresponding confidence interval determinedfrom a conventional estimation of a population proportion.
 18. Thenon-transitory computer readable media of claim 13 wherein the Bayesianpoint estimate of a population proportion has a smaller margin of errorfor any sample size or requires a smaller sample size for any desiredmargin of error than a conventionally estimated population proportion.19. The non-transitory computer readable media of claim 13 wherein amean of the population proportions and a mean of the sample proportionsover the plurality of options are each equal to an inverse of the numberof the plurality of options.
 20. The non-transitory computer readablemedia of claim 13 wherein the square of the correlation between thepopulation proportions and the sample proportions over the plurality ofoptions functions as a shrinkage coefficient.